Abstract
Using a quantitative methodology designed specifically for emerging economies, we measure the components of India’s economic growth over the period 1960–2005. Our approach accounts for time-varying parameters, transitional dynamics and non-linear trends. We find that increased productivity in the service sector, facilitated by a structural shift toward services, is the principal driver of India’s economic growth. Our measures also suggest that the allocation of inputs across sectors has not improved over this period, and in the case of labor appears to have significantly worsened. We further find that fluctuations in output around its trend are due primarily to fluctuations in sector-specific total factor productivity, with fluctuations in labor market distortions and labor taxes also playing important roles. In the period 1960–1980, productivity fluctuations in the agricultural sector are the dominant source of cycles. Since then, productivity fluctuations in the manufacturing and service sectors have been more important.
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Notes
Although sectoral shifts are a feature of developed economies as well, they are considered central to the process of development.
In the approach most similar to ours, Lahiri and Yi (2006) find non-linear transition paths for a deterministic, constant-parameter economy. Verma (2012) solves a constant-parameter model that generates sectoral shifts through unbalanced growth; her analysis, however, considers only the effects of productivity growth.
We model capital and labor frictions in different ways only to facilitate our derivations; in the end the frictions’ algebraic (as well as economic) effects are completely symmetric.
To facilitate our analysis of the model’s trends, our treatment of capital taxes—taxes are levied on capital itself—differs from the canonical BCA approach (Chari et al. 2007), where taxes are levied on investment.
Any effects that government spending might have on utility or production will be captured by correlations between government spending and the other wedges. Chari et al. (2007) provide several useful examples.
Bosworth et al.’s (2007) factor shares for agriculture include a component for land (0.25), which we divided evenly between capital and labor.
The variables \(\left\{ y_{t}^{*},\ell _{t}^{*}\right\} _{t}\) can be expressed as functions of \(\left\{ c_{t}^{*},k_{t}^{*}\right\} _{t}\) and the model’s forcing processes.
The only wrinkle is that with endogenous labor supply, the choice of \(c_{t+1} \) in Eq. (13) also affects \(y_{t+1}\), so that even when \(c_{t}\), \(k_{t}\), and \(k_{t+1}\) are known, there is no closed form solution for \(c_{t+1}\). However, it follows from Eq. (15) that \(y_{t+1}\) is decreasing in \(c_{t+1}\), so that given \(k_{t+1}\), the right-hand-side of Eq. (13) is monotonically decreasing in \(c_{t+1}\), and a numerical search is straightforward.
Allowing time variation introduces a small amount of imprecision into our solution; “Solving time-varying linear expectational difference equations” of Appendix provides a detailed discussion. Ignoring time variation, however, would arguably introduce more inaccuracy. For example, the parameter \(\theta \), the share of services, rises 27 percentage points over the sample period. In contrast, the magnitude of the solution errors appears to be less than 1 % of the consumption deviations.
In general, we assume that the trends follow our estimated trend equations until 2035, and then stay stable. The two exceptions are fiscal policies and depreciation, for which we did not feel comfortable making extended projections. We simply assume these variables stay at their 2005 (or 2006) trend values for the foreseeable future. “Estimated and projected trends” of Appendix shows the projected trends. Our results do not appear sensitive to these assumptions.
We are assuming that the productivity trends shifted slowly over the course of 6 years, rather than in a 1-period break. In all other respects, however, we treat the trend break as known in advance; our ability to divide the data into trend and deviations relies on this assumption. In contrast, Aguiar and Gopinath (2007) argue that trend breaks are the primary source of business cycles in emerging economies.
Bosworth et al. (2007, p. 39) find the amount of growth in India’s service sector productivity to be “quite puzzling”, and perhaps exaggerated by measurement error.
It bears noting that because the wedge series are not orthogonal, these experiments do not produce a true variance decomposition.
Verma (2012) considers this issue, and concludes (p. 173) “an export-led growth hypothesis of service sector growth is difficult to support.”
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Appendix: Background calculations
Appendix: Background calculations
1.1 Input prices
Assuming interiority, the first-order conditions for intermediate goods producers are
Under perfect competition and constant returns, the average cost of the final good will equal its price, and it follows from Eqs. (3a) to (3c) that
Combining these results with Eqs. (3a) to (3c) produces Eqs. (4a–5c).
1.2 Input aggregation
Combining Eqs. (1a–2), and inserting Eqs. (5a–5c) produces
and we can rewrite Eq. (2) as:
where
Combining Eqs. (4a) to (4c) with Eq. (7) produces
Letting \(\ell \) denote the average labor market “distortion”,
we can rewrite the preceding equations as
Inserting these results into Eq. (27) and rearranging to find Eq. (8):
with the term A defined as in the main text.
Combining Eqs. (4a), (7) and (28), we can rewrite the labor-leisure allocation condition as
This is Eq. (9) in the main text.
Combining Eqs. (3a) and (26) produces
Combining Eqs. (6) and (29) produces Eq. (11) in the main text.
1.3 Normalized production
Let lower case variables denote upper case variables divided by population and trend productivity, with \(c_{t}=C_{t}/\left( A^{*}_{t}N_{t}\right) \), and so on. The one exception is labor hours, where \(l_{t}=L_{t}/N_{t}\). Inserting these definitions into Eqs. (8) and (9), we get equation (15) in the main text:
with \(a_{t}\) and \(\phi _{t}\) defined as in the main text.
1.4 Linearization
Let hats (“\(\widehat{\; \;}\)”) denote deviations around the transition path. The tax rates, the allocation wedges (the \(\kappa \)’s and the \(\ell \)’s), the production parameters (\(\zeta \) and \(\phi \)), and the depreciation rate (\(\delta \)) are expressed as level deviations; in these cases \(\widehat{z}_{t}=z_{t}-z_{t}^{*}\). All other variables are expressed as log deviations; in these cases, \(\widehat{z}_{t}=\ln \left( z_{t}/z_{t}^{*}\right) \).
Consider the Euler equation
We can rewrite this expression as
where \(X_{A,t+1}^{*}=A_{t+1}^{*}/A_{t}^{*}\). Logging both sides, and assuming the deviations are small, one gets
Implicitly differentiating around trend values (“stars”, with “hats” set equal to zero), and noting that \(1/\left[ 1+\zeta _{t+1}(1+\kappa _{t+1}^{*})\lambda _{1,t+1}-\delta _{t+1}\right] =\lambda _{2,t+1}/(1+\tau _{kt}^{*})\), we get
Following Eq. (23), we replace \(\widehat{\tau }_{kt}\) with \(\rho _{k}\widehat{\widetilde{\tau }}_{k,t-1}\).
Next, consider the capital accumulation equation
which can be rewritten as
Implicit differentiation yields
or
To fill out these two difference equations, we substitute for output, using
Linearizing this equation requires us to consider the effects of the exponent deviations \(\widehat{\phi }_{t}\) and \(\widehat{\zeta }_{t}\). To see how this works, consider another expression for output:
This equality can be rewritten as
and taking logs yields
as \(\ln (a_{t}^{*})=0\).
To apply this approach to Eq. (30), we log both sides and implicitly differentiate:
Next, we substitute for the components of \(\widehat{a}_{t}\), \(\widehat{\kappa }_{t}, \widehat{\ell }_{t}, \widehat{\zeta }_{t}\) and \(\widehat{\phi }_{t}\). Consider first the expression for total factor productivity, \(\widehat{a}_{t}\). It follows from Eq. (10) that
Taking logs yields
or
Continuing, it follows from the definition of \(\Omega _{t}\) that:
Taking logs, and then implicitly differentiating, yields
Similarly,
Finally, it follows from Eqs. (10) and (18) that
so that
Next, we consider the sectoral distortions, \(\kappa _{t}\) and \(\ell _{t}\). Recall the capital distortion:
Implicit differentiation yields
Similarly
Finally, we express the composite parameters \(\zeta _{t}\) and \(\phi _{t}\) as functions of the share parameters \(\psi _{t}\), \(\eta _{t}\) and \(\theta _{t}\). Recall that
yielding
1.5 Solving time-varying linear expectational difference equations
Our approach most closely follows that of Klein (2000) (and to a lesser extent Sims 2002), although we use the eigenvalue-eigenvector decomposition introduced by Blanchard and Kahn (1980) (as well as their notation), rather than the generalized Schur decomposition. We also incorporate elements of the MDS approaches used by Broze et al. (1985), Farmer (1993) and Farmer and Guo (1994), as well as insights from King et al. (2002).
1.5.1 The basic solution
Consider the following system:
where \({\mathbf {v}}_{t}\) is an \((n\times 1)\) vector, \({\mathbf {A}}_{t}\) is an \((n\times n)\) time-dependent matrix, and \(E_{t}\left( {\mathbf {.}}\right) \) is the usual conditional expectations operator. The vector \({\mathbf {v}}_{t}\) can be decomposed into the \((n_{1}\times 1)\) control vector \({\mathbf {x}}_{t}\) and the \((n_{2}\times 1)\) state vector \({\mathbf {p}}_{t}\). In particular, \({\mathbf {p}}_{t}\) is restricted by
where \(\left\{ {\mathbf {d}}_{t+1}\right\} _{t=0}^{\infty }\) is a covariance stationary Martingale difference sequence. In contrast, \({\mathbf {x}}_{t}\) needs only to obey some standard boundedness conditions; \({\mathbf {g}}_{t+1}\equiv {\mathbf {x}}_{t+1}-E_{t}\left( {\mathbf {x}}_{t+1}\right) \) is otherwise unrestricted. Using these definitions, we can rewrite Eq. (31) as
subject to the restrictions in Eq. (32). In our case, the control vector \({\mathbf {x}}_{t}\) has one variable, the consumption deviation (\(\widehat{c}_{t}\)), so that \(n_{1}=1\). The remaining variables, the capital and wedge deviations, are elements of \({\mathbf {p}}_{t}\).
The next step is to diagonalize \({\mathbf {A}}_{t}\):
where: the matrix \({\mathbf {B}}_{t}\) contains the eigenvectors of \({\mathbf {A}}_{t}\); the matrix \({\mathbf {J}}_{t}\) is a diagonal matrix holding the associated eigenvalues; and \({\mathbf {C}}_{t}={\mathbf {B}}_{t}^{-1}\). (In our case, a simple diagonalization always works). Assume that the eigenvalues are sorted by size in descending order, and let \(m_{1}\) denote the number of eigenvalues of magnitude greater than 1. In the standard saddle-path case, \(m_{1}=n_{1}\). We will hold this assumption throughout our analysis; readers interested in other configurations can consult the references listed above. With saddle-path stability, we can partition \({\mathbf {B}}_{t}\), \({\mathbf {J}}_{t}\), and \({\mathbf {C}}_{t},\) as:
Premultiplying Eq. (33) by \({\mathbf {C}}_{t}\) yields the transformed system
Because the timing of the transformation is important, we use tildes to denote transformed variables with time “mismatches”.
Because the elements of \({\mathbf {J}}_{1t}\) are bigger than 1 in magnitude, the non-explosive solution to the first row of Eq. (34) is to set
It immediately follows that
But because \({\mathbf {d}}_{t+1}\) is given, it must be the case that
and
where \({\mathbf {I}}_{n_{2}}\) is an identity matrix of size \(n_{2}\).
1.5.2 The effects of time variation
Equation (35) implies that the innovation to the control variable \({\mathbf {x}}_{t}\) is a linear function of the innovations to the state variable \({\mathbf {p}}_{t}\). The same logic, however, applies to the variables \({\mathbf {x}}_{t}\) and \({\mathbf {p}}_{t}\) themselves. The fact that \({\mathbf {p}}_{t}\) is pre-determined at time t, along with the non-explosiveness restriction \({\mathbf {y}}_{t}={\mathbf {0}}\) , implies that
Comparing Eqs. (35) and (36) reveals a timing inconsistency: time-t innovations are “stabilized” using time-\(t-1\) coefficients, while the variables themselves are stabilized using time-t coefficients. To see how this plays out, consider the system:
Suppose further that: \({\mathbf {v}}_{0}=\) \({\mathbf {0}}\); \({\mathbf {d}}_{t}={\mathbf {0}}\), \(\forall t\ne 1\); and \({\mathbf {d}}_{1}\ne {\mathbf {0}}\). This yields:
But it should also be the case that
Following Klein (2000), we can show that Eq. (36) generates a bounded solution. In particular,
Moreover,
As a result,
because, as noted by Klein (2000, p. 1418), \({\mathbf {C}}_{t}{\mathbf {B}}_{t}={\mathbf {I}}\).
Note that
where the “stabilized” transition matrix \({\mathbf {A}}_{t}^{*}\) has been purged of its explosive eigenvalues:
In short, applying Eq. (36) is equivalent to updating Eq. (33) with a non-explosive transition matrix. This result does not hold if we use \({\mathbf {H}}_{t-1}\) from Eq. (35), as \({\mathbf {A}}_{t}\) contains \({\mathbf {C}}_{1t}\), while \({\mathbf {H}}_{t-1}\) contains \({\mathbf {B}}_{2,t-1}\). On the other hand, using Eq. (35) bests captures the transition dynamics in effect at time t. Our solution is this:
-
1.
Given \({\mathbf {p}}_{t}\), use Eq. (36) to find \({\mathbf {x}} _{t}\).
-
2.
Given \(\left( {\mathbf {x}}_{t}^{\prime },{\mathbf {p}}_{t}^{\prime }\right) ^{\prime }\), use the bottom \(n_{2}\) rows of Eqs. (33) or (37) to find \({\mathbf {p}}_{t+1}\). Return to step 1.
Using this approach means that \({\mathbf {x}}_{t+1}\) is not entirely consistent with the dynamics implied by Eqs. (33) or (37). In our context, this means that consumption does not perfectly satisfy the linearized Euler equation. The error appears to be less than 1 % of consumption, however, which is small relative to some observed parameter changes.
1.6 Estimated and projected trends
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Jones, J.B., Sahu, S. Transition accounting for India in a multi-sector dynamic general equilibrium model. Econ Change Restruct 50, 299–339 (2017). https://doi.org/10.1007/s10644-016-9190-1
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DOI: https://doi.org/10.1007/s10644-016-9190-1